# Computing and visualizing LDA in R

As I have described before, Linear Discriminant Analysis (LDA) can be seen from two different angles. The first classify a given sample of predictors ${x}$ to the class ${C_l}$ with highest posterior probability ${\pi(y = C_l|x)}$. It minimizes the total probability of misclassification. To compute ${\pi(y = C_l|x)}$ it uses Bayes’ rule and assume that ${\pi(x|y = C_l)}$ follows a Gaussian distribution with class-specific mean ${\mu_l}$ and common covariance matrix ${\Sigma}$. The second tries to find a linear combination of the predictors that gives maximum separation between the centers of the data while at the same time minimizing the variation within each group of data.

The second approach [1] is usually preferred in practice due to its dimension-reduction property and is implemented in many R packages, as in the lda function of the MASS package for example. In what follows, I will show how to use the lda function and visually illustrate the difference between Principal Component Analysis (PCA) and LDA when applied to the same dataset.

Using lda from MASS R package

As usual, we are going to illustrate lda using the iris dataset. The data contains four continuous variables which correspond to physical measures of flowers and a categorical variable describing the flowers’ species.

require(MASS)

data(iris)

Sepal.Length Sepal.Width Petal.Length Petal.Width Species
1          5.1         3.5          1.4         0.2  setosa
2          4.9         3.0          1.4         0.2  setosa
3          4.7         3.2          1.3         0.2  setosa


An usual call to lda contains formula, data and prior arguments [2].

r <- lda(formula = Species ~ .,
data = iris,
prior = c(1,1,1)/3)


The . in the formula argument means that we use all the remaining variables in data as covariates. The prior argument sets the prior probabilities of class membership. If unspecified, the class proportions for the training set are used. If present, the probabilities should be specified in the order of the factor levels.

> r$prior setosa versicolor virginica 0.3333333 0.3333333 0.3333333 > r$counts
setosa versicolor  virginica
50         50         50

> r$means Sepal.Length Sepal.Width Petal.Length Petal.Width setosa 5.006 3.428 1.462 0.246 versicolor 5.936 2.770 4.260 1.326 virginica 6.588 2.974 5.552 2.026 > r$scaling
LD1         LD2
Sepal.Length  0.8293776  0.02410215
Sepal.Width   1.5344731  2.16452123
Petal.Length -2.2012117 -0.93192121
Petal.Width  -2.8104603  2.83918785

> r$svd [1] 48.642644 4.579983  As we can see above, a call to lda returns the prior probability of each class, the counts for each class in the data, the class-specific means for each covariate, the linear combination coefficients (scaling) for each linear discriminant (remember that in this case with 3 classes we have at most two linear discriminants) and the singular values (svd) that gives the ratio of the between- and within-group standard deviations on the linear discriminant variables. prop = r$svd^2/sum(r$svd^2) > prop [1] 0.991212605 0.008787395  We can use the singular values to compute the amount of the between-group variance that is explained by each linear discriminant. In our example we see that the first linear discriminant explains more than ${99\%}$ of the between-group variance in the iris dataset. If we call lda with CV = TRUE it uses a leave-one-out cross-validation and returns a named list with components: • class: the Maximum a Posteriori Probability (MAP) classification (a factor) • posterior: posterior probabilities for the classes. r2 <- lda(formula = Species ~ ., data = iris, prior = c(1,1,1)/3, CV = TRUE) > head(r2$class)
[1] setosa setosa setosa setosa setosa setosa
Levels: setosa versicolor virginica

> head(r2$posterior, 3) setosa versicolor virginica 1 1 5.087494e-22 4.385241e-42 2 1 9.588256e-18 8.888069e-37 3 1 1.983745e-19 8.606982e-39  There is also a predict method implemented for lda objects. It returns the classification and the posterior probabilities of the new data based on the Linear Discriminant model. Below, I use half of the dataset to train the model and the other half is used for predictions. train <- sample(1:150, 75) r3 <- lda(Species ~ ., # training model iris, prior = c(1,1,1)/3, subset = train) plda = predict(object = r, # predictions newdata = iris[-train, ]) > head(plda$class) # classification result
[1] setosa setosa setosa setosa setosa setosa
Levels: setosa versicolor virginica

> head(plda$posterior, 3) # posterior prob. setosa versicolor virginica 3 1 1.463849e-19 4.675932e-39 4 1 1.268536e-16 3.566610e-35 5 1 1.637387e-22 1.082605e-42 > head(plda$x, 3) # LD projections
LD1        LD2
3 7.489828 -0.2653845
4 6.813201 -0.6706311
5 8.132309  0.5144625


Visualizing the difference between PCA and LDA

As I have mentioned at the end of my post about Reduced-rank DA, PCA is an unsupervised learning technique (don’t use class information) while LDA is a supervised technique (uses class information), but both provide the possibility of dimensionality reduction, which is very useful for visualization. Therefore we would expect (by definition) LDA to provide better data separation when compared to PCA, and this is exactly what we see at the Figure below when both LDA (upper panel) and PCA (lower panel) are applied to the iris dataset. The code to generate this Figure is available on github.

Although we can see that this is an easy dataset to work with, it allow us to clearly see that the versicolor specie is well separated from the virginica one in the upper panel while there is still some overlap between them in the lower panel. This kind of difference is to be expected since PCA tries to retain most of the variability in the data while LDA tries to retain most of the between-class variance in the data. Note also that in this example the first LD explains more than ${99\%}$ of the between-group variance in the data while the first PC explains ${73\%}$ of the total variability in the data.

Closing remarks

Although I have not applied it on my illustrative example above, pre-processing [3] of the data is important for the application of LDA. Users should transform, center and scale the data prior to the application of LDA. It is also useful to remove near-zero variance predictors (almost constant predictors across units). Given that we need to invert the covariance matrix, it is necessary to have less predictors than samples. Attention is therefore needed when using cross-validation.

References:

[1] Venables, W. N. and Ripley, B. D. (2002). Modern applied statistics with S. Springer.
[2] lda (MASS) help file.
[3] Kuhn, M. and Johnson, K. (2013). Applied Predictive Modeling. Springer.

# Reduced-rank discriminant analysis

In a previous post, I described linear discriminant analysis (LDA) using the following Bayes’ rule

$\displaystyle \pi(y=C_l | x) = \frac{\pi(y=C_l) \pi(x|y=C_l)}{\sum _{l=1}^{C}\pi(y=C_l) \pi(x|y=C_l)}$

where ${\pi(y=C_l)}$ is the prior probability of membership in class ${C_l}$ and the distribution of the predictors for a given class ${C_l}$, ${\pi(x|y=C_l)}$, is a multivariate Gaussian, ${N(\mu_l, \Sigma)}$, with class-specific mean ${\mu_l}$ and common covariance matrix ${\Sigma}$. We then assign a new sample ${x}$ to the class ${C_k}$ with the highest posterior probability ${\pi(y=C_k | x)}$. This approach minimizes the total probability of misclassification [1].

Fisher [2] had a different approach. He proposed to find a linear combination of the predictors such that the between-class covariance is maximized relative to the within-class covariance. That is [3], he wanted a linear combination of the predictors that gave maximum separation between the centers of the data while at the same time minimizing the variation within each group of data.

Variance decomposition

Denote by ${X}$ the ${n \times p}$ matrix with ${n}$ observations of the ${p}$-dimensional predictor. The covariance matrix ${\Sigma}$ of ${X}$ can be decomposed into the within-class covariance ${W}$ and the between-class covariance ${B}$, so that

$\displaystyle \Sigma = W + B$

where

$\displaystyle W = \frac{(X - GM)^T(X-GM)}{n-C}, \quad B = \frac{(GM - 1\bar{x})^T(GM - 1\bar{x})}{C-1}$

and ${G}$ is the ${n \times C}$ matrix of class indicator variables so that ${g_{ij} = 1}$ if and only if case ${i}$ is assigned to class ${j}$, ${M}$ is the ${C \times p}$ matrix of class means, ${1\bar{x}}$ is the ${n \times p}$ matrix where each row is formed by ${\bar{x}}$ and ${C}$ is the total number of classes.

Fisher’s approach

Using Fisher’s approach, we want to find the linear combination ${Z = a^TX}$ such that the between-class covariance is maximized relative to the within-class covariance. The between-class covariance of ${Z}$ is ${a^TBa}$ and the within-class covariance is ${a^TWa}$.

So, Fisher’s approach amounts to maximizing the Rayleigh quotient,

$\displaystyle \underset{a}{\text{max }} \frac{a^TBa}{a^T W a}$

[4] propose to sphere the data (see page 305 of [4] for information about sphering) so that the variables have the identity as their within-class covariance matrix. That way, the problem becomes to maximize ${a^TBa}$ subject to ${||a|| = 1}$. As we saw in the post about PCA, this is solved by taking ${a}$ to be the eigenvector of B corresponding to the largest eigenvalue. ${a}$ is unique up to a change of sign, unless there are multiple eigenvalues.

Dimensionality reduction

Similar to principal components, we can take further linear components corresponding to the next largest eigenvalues of ${B}$. There will be at most ${r = \text{min}(p, C-1)}$ positive eigenvalues. The corresponding transformed variables are called the linear discriminants.

Note that the eigenvalues are the proportions of the between-class variance explained by the linear combinations. So we can use this information to decide how many linear discriminants to use, just like we do with PCA. In classification problems, we can also decide how many linear discriminants to use by cross-validation.

Reduced-rank discriminant analysis can be useful to better visualize data. It might be that most of the between-class variance is explained by just a few linear discriminants, allowing us to have a meaningful plot in lower dimension. For example, in a three-class classification problem with many predictors, all the between-class variance of the predictors is captured by only two linear discriminants, which is much easier to visualize than the original high-dimensional space of the predictors.

Regarding interpretation, the magnitude of the coefficients ${a}$ can be used to understand the contribution of each predictor to the classification of samples and provide some understanding and interpretation about the underlying problem.

RRDA and PCA

It is important to emphasize the difference between PCA and Reduced-rank DA. The first computes linear combinations of the predictors that retain most of the variability in the data. It does not make use of the class information and is therefore an unsupervised technique. The latter computes linear combinations of the predictors that retain most of the between-class variance in the data. It uses the class information of the samples and therefore belongs to the group of supervised learning techniques.

References:

[1] Welch, B. L. (1939). Note on discriminant functions. Biometrika, 31(1/2), 218-220.
[2] Fisher, R. A. (1936). The use of multiple measurements in taxonomic problems. Annals of eugenics, 7(2), 179-188.
[3] Kuhn, M. and Johnson, K. (2013). Applied Predictive Modeling. Springer.
[4] Venables, W. N. and Ripley, B. D. (2002). Modern applied statistics with S. Springer.

# Discriminant Analysis

According to Bayes’ Theorem, the posterior distribution that a sample ${x}$ belong to class ${l}$ is given by

$\displaystyle \pi(y=C_l | x) = \frac{\pi(y=C_l) \pi(x|y=C_l)}{\sum _{l=1}^{C}\pi(y=C_l) \pi(x|y=C_l)} \ \ \ \ \ (1)$

where ${\pi(y = C_l)}$ is the prior probability of membership in class ${C_l}$ and ${\pi(x|y=C_l)}$ is the conditional probability of the predictors ${x}$ given that data comes from class ${C_l}$.

The rule that minimizes the total probability of misclassification says to classify ${x}$ into class ${C_k}$ if

$\displaystyle \pi(y = C_k)\pi(x|y=C_k) \ \ \ \ \ (2)$

has the largest value across all of the ${C}$ classes.

There are different types of Discriminant Analysis and they usually differ on the assumptions made about the conditional distribution ${\pi(x|y=C_l)}$.

Linear Discriminant Analysis (LDA)

If we assume that ${\pi(x|y=C_l)}$ in Eq. (1) follows a multivariate Gaussian distribution ${N(\mu_l, \Sigma)}$ with a class-specific mean vector ${\mu_l}$ and a common covariance matrix ${\Sigma}$ we have that the ${\log}$ of Eq. (2), referred here as discriminant function, is given by

$\displaystyle x^T\Sigma^{-1}\mu_l - 0.5\mu_l^T \Sigma ^{-1}\mu_l + \log (\pi(y = C_l)),$

which is a linear function in ${x}$ that defines separating class boundaries, hence the name LDA.

In practice [1], we estimate the prior probability ${\pi(y = C_l)}$, the class-specific mean ${\mu_l}$ and the covariance matrix ${\Sigma}$ by ${\hat{\pi}_l}$, ${\hat{\mu}_l}$ and ${\hat{\Sigma}}$, respectively, where:

• ${\hat{\pi}_l = N_l/N}$, where ${N_l}$ is the number of class ${l}$ observations and ${N}$ is the total number of observations.
• ${\hat{\mu}_l = \sum_{\{i:y_i=l\}} x_i/N_l}$
• ${\hat{\Sigma} = \sum_{l=1}^{C} \sum_{\{i:y_i=l\}} (x_i - \hat{\mu}_l)(x_i - \hat{\mu}_l)^T/(N-C)}$

If instead we assume that ${\pi(x|y=C_l)}$ in Eq. (1) follows a multivariate Gaussian ${N(\mu_l, \Sigma _l)}$ with class-specific mean vector and covariance matrix we have a quadratic discriminant function

$\displaystyle -0.5 \log |\Sigma _l| - 0.5(x - \mu_l)^T \Sigma _l ^{-1}(x - \mu_l) + \log (\pi(y = C_l)),$

and the decision boundary between each pair of classes ${k}$ and ${l}$ is now described by a quadratic equation.

Notice that we pay a price for this increased flexibility when compared to LDA. We now have to estimate one covariance matrix for each class, which means a significant increase in the number of parameters to be estimated. This implies that the number of predictors needs to be less than the number of cases within each class to ensure that the class-specific covariance matrix is not singular. In addition, if the majority of the predictors in the data are indicators for discrete categories, QDA will only be able to model these as linear functions, thus limiting the effectiveness of the model [2].

Regularized Discriminant Analysis

Friedman ([1], [3]) proposed a compromise between LDA and QDA, which allows one to shrink the separate covariances of QDA toward a common covariance as in LDA. The regularized covariance matrices have the form

$\displaystyle \Sigma (\alpha) = \alpha \Sigma _l + (1-\alpha) \Sigma, \ \ \ \ \ (3)$

where ${\Sigma}$ is the common covariance matrix as used in LDA and ${\Sigma _l}$ is the class-specific covariance matrix as used in QDA. ${\alpha}$ is a number between ${0}$ and ${1}$ that can be chosen based on the performance of the model on validation data, or by cross-validation.

It is also possible to allow ${\Sigma}$ to shrunk toward the spherical covariance

$\displaystyle \Sigma(\gamma) = \gamma \Sigma + (1 - \gamma) \sigma ^2 I,$

where ${I}$ is the identity matrix. The equation above means that, when ${\gamma = 0}$, the predictors are assumed independent and with common variance ${\sigma ^2}$. Replacing ${\Sigma}$ in Eq. (3) by ${\Sigma(\gamma)}$ leads to a more general family of covariances ${\Sigma(\alpha, \gamma)}$ indexed by a pair of parameters that again can be chosen based on the model performance on validation data.

References:

[1] Hastie, T., Tibshirani, R., Friedman, J. (2009). The elements of statistical learning: data mining, inference and prediction. Springer.
[2] Kuhn, M. and Johnson, K. (2013). Applied Predictive Modeling. Springer.
[3] Friedman, J. H. (1989). Regularized discriminant analysis. Journal of the American statistical association, 84(405), 165-175.